1. No category

An assessment of sample-based estimators of tree species
richness in two wet tropical forest compartments in Panama
and India
S. MAGNUSSEN, R. PÉLISSIER, F. HE AND B.R. RAMESH
Natural Resources Canada, Canadian Forest Service, 506 West Burnside Rd., Victoria
BC V8Z 1M5, CANADA.
© 2006, Her Majesty the Queen in Right of Canada
Email: [email protected]
Note: This is a pdf preprint of the final submission made to the journal before
publication. There may be minor editorial differences between this version and the final
version published in International Forestry Review 8(4): 417-431.
The final published version of this article can be viewed on the journal web site at
http://www.atypon-link.com/CFA/doi/abs/10.1505/ifor.8.4.417
1
SUMMARY
We assessed the performance of ten incidence-based estimators of tree species
richness in simulated simple random sampling with fixed-area plots. Stem diameterlimited tree species and location data came from two species-rich wet tropical forest
compartments in Panama and India. Lower limits of stem diameter were 1 cm and 30 cm,
respectively. Estimators varied widely in their estimates of richness and their rankings
changed frequently across sites and sample designs. A gamma-Poisson estimator was
overall best according to a performance score of three accuracy statistics and sample size.
However, until corroborated by further studies Chao’s 1981 non-parametric estimator is
recommended for forest inventories with fixed-area plots.
Keywords: probability of occurrence, variance estimator, sample coverage, sample size,
plot size, mixed-models.
2
INTRODUCTION
An important indicator of sustainable forest development is the conservation of
biological diversity (McDonald and Lane 2004). The number of tree species (richness) in
our forests is a key component of this diversity. Obtaining an unbiased and precise
estimate of the number of forest tree species currently growing in a region, state or
country poses a challenge. Species lists compiled from historic data, anecdotal evidence
or tree distribution maps may not reflect current reality (Guralnick and Van Cleve 2005).
It is, of course, the rare tree species that pose the challenge. A direct monitoring or survey
of the status of all rare species in each area of interest would provide the needed data but
this option is rarely realistic (Acharya, Bhattarai, de Gier et al. 2000, Gimaret-Carpentier,
Pelissier, Pascal et al. 1998, Green 1993, Venette, Moon, and Hutchison 2002, Yatracos
1995).
Existing forest surveys would ideally provide an estimate of the number of tree
species in a population of interest. A long history of research on estimation of the number
of species in an area (Arrhenius 1921, Evans, Clark, and Brand 1955, Fisher, Corbet, and
Williams 1943) has provided us with a plethora of estimators and estimation procedures
(for examples, Chao and Bunge 2002, Chazdon, Colwell, Denslow et al. 1998, Dorazio
and Royle 2003, Skov and Lawesson 2000, Tackaberry, Brokaw, Kellman et al. 1997,
and Walther and Morand 1998). Each estimator rests on a set of assumptions about the
population and the sampling protocol (Bunge and Fitzpatrick 1993). It is a common
observation that estimates obtained from these estimators are sensitive to both the
structure of the sampled population and the sample design (Brose, Martinez, and
Williams 2003, Colwell, Mao, and Chang 2004, Keating, Quinn, Ivie et al. 1998). Rare
species, easily missed in a survey, exert a disproportionate influence on the results (Link
2003, Mao and Colwell 2005). Samples with a poor representation of rare species cannot
be expected to yield reliable estimates of richness. It is generally recognized that the
estimation problem is intransigent and that estimators are biased (Bunge and Fitzpatrick
1993).
Can we expect a forest survey - designed to provide accurate estimates of the area
of different forest cover-type classes, wood volume, and biomass – to provide a reliable
estimate of the number of tree species in a forest? Apparently the answer will depend
both on the forest and the sample strategy (sampling design and estimation). Experience
with sample-based estimation of tree species richness is limited. Schreuder, Williams,
3
and Reich (1999) assessed 10 modifications of Chao and Lee’s non-parametric
estimators by sampling two large data sets with 4060 forest inventory plots from
Missouri and 12260 from Minnesota, respectively. Sample sizes in the order of 500 to
700 were deemed necessary to keep bias below 15%. Sample sizes of 80 produced a
(negative) bias of about 40%. Palmer (1990 and 1991) investigated eight estimators in
sampling (trees, forbs, and herbs) from 30 20 m × 20 m plots in the Duke Forest (North
Carolina, USA) with 40 circular 2 m2 samples taken from each plot. The second-order
jackknifed and the bootstrap (Smith and van Belle 1984) performed best in terms of
accuracy and precision. Hellmann and Fowler (1999) simulated sampling with 5 m × 5 m
plots within five forested 0.4 ha plots in Michigan. Bias of estimates obtained with
jackknifed and bootstrap estimators depended strongly on sample size with a switch from
negative to positive bias as the sampled area surpassed approximately 35% of the total
area. The second-order jackknifed estimator was the best for low-intensity sampling (<
10% of area sampled). Gimaret-Carpentier, Pélissier, Pascal et al. (1998) compared the
behavior of Chaos’ and the generalized jackknifed estimators of richness in two wet
tropical forests under random and systematic sampling with cluster-sizes of 1, 10, 50, and
100 trees. Chao’s estimator(s) were superior to the generalized jackknifed estimator(s)
and systematic sampling was more efficient than random sampling. Cluster-size effects
were restricted to designs with less than a total of 400 sampled trees. Krishnamani,
Kumar, and Harte (2004) reached a realistic estimate of 893 for the number of tree
species in 60 000 km2 of the forests in the Western Ghats (India) from just 48 (0.25 ha)
plots. An index of species similarity between a pair of plots was cast as a non-linear
function of inter-plot distances and used in a plug-in formula for the species-area
relationship. Excellent summaries of techniques and methods for assessing species
richness have been given by, for examples, Chazdon, Colwell, Denslow et al. (1998),
Condit, Hubbell, Lafrankie et al. (1996) and Tackaberry, Brokaw, Kellman et al. (1997).
It is well known that only an intensive survey of a population can generate an accurate
estimate of richness (for example Gimaret-Carpentier, Pélissier, Pascal et al. 1998).
The objective of this study is to assess the performance (bias, precision, and
accuracy) of six non-parametric and four model-based richness estimators in the context
of a forest inventory. Based on the outcome of this assessment we discuss the pros and
cons of attempting to estimate tree species richness from an inventory sample, and we
make a recommendation for those who decide to produce an estimate of forest tree
species richness from a forest inventory sample.
4
A forest inventory typically samples only a small fraction of a forest. Sample
sizes are sufficient to estimate the mean or total of a quantitative trait with a desired
accuracy and precision. It is at these low sampling intensities that estimates of richness
depends most on the chosen estimator and estimators are needed the most due to a
considerable negative bias in the observed richness (Walther and Moore 2005). Our
assessment of the estimators is based on simulated simple random sampling with fixedarea plots in two tree species rich wet tropical forest compartments with a high frequency
of rare tree species. Fixed-area plots are commonplace in forest inventories (Köhl,
Magnussen, and Marchetti 2006). Simple random sampling with fixed-area plots affords
unbiased estimates of stem inclusion probabilities. All information needed for an
estimation of tree species richness are contained in the species incidence statistic (the
number of sample plots containing a given species). Consequently only incidence-based
estimators are considered (Fattorini in press, Hurlbert 1971). Estimators tailored towards
a comparison of independent estimates of species richness (rarefaction) or temporal
trends are beyond the scope of this study. We also screened out estimators that are
suitable only for large sample sizes, notably estimators based on species-area and
species-accumulation curves (for example, Ugland, Gray, and Ellingsen 2003, and
Dorazio and Royle 2005 for a further discussion of this point).
Our final selection of incidence-based estimators of species richness has been
based on a recent review (Walther and Moore 2005) and experience from a preliminary
study. We anticipate that an estimator that does well on two species rich sites will also do
well on sites with fewer tree species, although species richness can influence estimator
performance (Walther and Morand 1998).
A good estimator is one that across a range of sample sizes and a range of
populations consistently produces estimates that are closest to the actual richness
(Schreuder, Williams, and Reich 1999, Walther and Morand 2005). In our assessment we
take the position that an estimator should have a low risk of producing and inflated
estimate of richness. This position is based on a mix of statistical and practical
implications of an inflated estimate. Not only does the standard error of a richness
estimate increase in proportion to the estimate of the unseen number of species, but the
reliability of this estimates declines at an exponential rate (Mao and Colwell 2005).
Secondly, the credibility of a forest inventory agency can be irrevocably damaged if it
produces an estimate that later has to be retracted as more information becomes available.
5
MATERIAL AND METHODS
Data
Two stem-mapped stand-level data sets will be used for the assessment of samplebased estimators of species richness. The first is from the Kadamakal Reserve Forest
(Kadagu District, Karnatiaka State, India) near the village of Uppangala in the Western
Ghats mountain range (12º 30' N, 75º 39' W; 500-600 m altitude). The forest type is
Dipterocarpus indicus-Kingiodendron pinnatum-Humboldtia brunonis (Pascal 1982).
Within a 28 ha forest compartment five 20 m wide north-south oriented strips 100 m
apart and 180 to 370 m long were inventoried (Pascal and Pélissier 1996). The species
and the spatial location were determined for all trees with a diameter at breast height
larger than 30 cm. In the inventoried area of 3.12 ha Pascal and Pélissier (1996) found
1981 such trees (635 trees per ha) with a basal area of 39.7 m2 ha-1. Ninety-three species
belonging to 31 families were identified in the five strips. An additional 12 species were
seen in the 28 ha stand but not in the five strips. Figure 1 lends an impression of the
species distribution in the five strips. Here the relative number of species is plotted
against the relative number of 5 m × 20 m plots in which they occur. The distribution is
typical of species rich wet tropical forests. Forty-six percent of the species were found in
less than 1% of the plots and 90% in less than 13%. On average a species was found in
just 5% of the plots. One species was seen in 62% of the plots. Pascal and Pélissier
(1996) estimated Simpson’s diversity index at 0.92 (i.e. about 92 pairs of trees selected at
random out of a 100 are composed of different species) and that of Shannon’s at 4.56
(compare to a maximum value of 6.54). The number of different species in a 100 m2 plot
varied from a low of 0 in two empty plots to a high of 13 with a mean and median of 6
(Figure 2). We refer to this site as WGHAT in the results and the discussion.
[FIGURES 1 AND 2 HERE]
The second data set is from a rich old-growth stem-mapped wet tropical forest
compartment dominated by Leguminosae and Bomabcaceae (area 1000 m × 500 m = 50
ha) on the Barro Colorado Island in the Panama Canal (Condit, Hubbell, Lafrankie et al.
1996, He and Hubbell 2003). Data from the 1990 census are used in this study. A total of
220 000 trees (4400 ha-1) of all sizes representing 301 species were identified. Further
details are in He and Hubbell (2003). In Figure 3 the relative number of species is plotted
against the relative number of square 156-m2 plots in which they occur. The distribution
is similar to that of WGHAT and also typical of species rich wet tropical forests. Thirtysix percent of the species were found in less than 1% of the plots, 50% in less than 2%,
6
and 90% in less than 30% of the plots. One species was found in all plots. The number of
different species in a 156 m2 plot varied from a low of 5 to a high of 52 with a mean and
median of 30 (Figure 4). We shall refer to this site as BCI.
[FIGURES 3 AND 4 HERE]
Sampling designs and sample statistics
In WGHAT the five 20 m wide survey lines totaling 1560 m in length were
subdivided into 312 100 m2 rectangular (5 m × 20 m) plots. Simple random sampling
with sample sizes n = 10, 15,..., 30 plots without replacement was simulated. Accordingly
between 3.2% and 9.6% of the area was sampled. In BCI the 50 ha area was tessellated
into 3200, 1250, and 800 square plots with side lengths of 12.5 m, 20 m, and 25 m,
respectively. Simple random sampling with n = 40, 60,..., 140 plots without replacement
was simulated.
Sampling followed by estimation of tree species richness (S) was repeated 2000
times for each combination of sample and plot size. Estimates of variance were corrected
for finite population size by a factor 1-fpc with f pc equal to the proportion of the area
sampled (Cochran, 1977).
Let SOBS be the number of species encountered in n sample plots. Encountered
species are labeled by an index i ( i = 1,..., SOBS ) . The sample data consist of a size
SOBS × n binary matrix δ with element δ ij = 1 if the ith species occurred in the jth plot
and zero otherwise. The number of plots containing the ith species is δ i = ∑ j =1 δ ij and
n
the proportion of sample plots with this species is pi . For incidence-based estimation of
richness the incidence vector f = ( f1 ,K, f n ) with f k = # {δ i = k} contains all the needed
information. We have SOBS = ∑ k =1 f k . Simply put, f is indexing the number of species
n
encountered in 1, 2,..., n plots (Esty 1982, Starr 1979). Species not encountered in any
sampling plot are captured by f0, the number of ‘missed’ species. We have S = SOBS + f 0 .
7
The total number of times one of the observed species were encountered in the n
plots is referred to as ncatch , in reference to the ‘total catch’, in animal capture-recapture
parlance (Burnham and Overton 1978, Chao 1989). ncatch is a design-dependent random
variable that also depends on the population structure. We have
S
n
nˆcatch = ∑ i =OBS
δ i =∑ k =1 k × f k . The proportion of the ith species in ncatch is
1
πˆi =
1
∑δ
n
∑
SOBS
πˆi ≡ 1 . We shall denote π i as the design and population
nˆcatch
dependent ‘catch’ probability of the ith species. A design-unbiased estimator of the
sampling variance of SOBS is not available. The distribution of SOBS has been assumed
2
−1
as an estimator
Poisson with a mean and a variance equal to SOBS. We propose SOBS
× ncatch
of the sampling variance on the grounds that ncatch / SOBS is the average number of plots
per unique species in the sample.
j
ij
with
i =1
Estimators of richness
Nine incidence-based estimators of richness are evaluated. Only a sketch of the
estimators is given here. Details are found in the provided references. A software
program for the estimators has been written in MATHEMATICA® (Wolfram Research
2005) and is available upon request to the senior author.
Petersen’s capture-recapture estimator (PET)
Petersen’s capture-recapture estimator (see Thompson, 1992 page 214 EQ 3) of
richness is
(1)
(2)
⎡
SOBS
(1) ⎤
ˆ
S PET = E ⎢ η × (1)∩(2) × SOBS
⎥
⎣ SOBS
⎦
(1)
where SOBS
denotes the number of species found only in the first-half of a random split of
(2)
the sample plots, SOBS
is the number of species found only in the second half of the
8
(1) ∩ (2)
random split, SOBS
is the number of species found in both halves,
(1)
(2)
(1)
, and E stands for the expectation over random splits (here 1000).
η = ( SOBS
+ SOBS
) / SOBS
The bootstrap estimator (BOOT)
Smith and van Belle (1984) were the first to suggest a bootstrap estimation of
richness. A bootstrap sample of size n is drawn with replacement from the sample
r
records. Let S BOOT
be the number of unique species in the rth bootstrap sample. The
r
− SOBS ⎤⎦ over all possible bootstrap samples is an estimate of
expected difference E ⎡⎣ S BOOT
the bias in the observed richness, and when added to the observed richness it yields the
bootstrap estimate of richness. Smith and van Belle (1984) also detail the estimator we
use for the sampling variance of the bootstrap estimate.
The generalized jackknifed estimator (JKk)
The generalized jackknifed estimator of richness is a linear combination of
conventional jackknifed estimators (Sharot 1976). The order of a generalized jackknifed
estimator (k) defines the linear combination of conventional jackknifed estimators that
enters into the generalization. The estimator is
(2)
1 k
j ⎛k ⎞
k
SˆJKk = ∑ ( −1) ⎜ ⎟ ( n − j ) Sˆ− j
k ! j =0
⎝ j⎠
(
)
where Sˆ− j is the conventional leave j-out jackknifed estimator of richness Sˆ0 ≡ SOBS .
Burnham and Overton (1978) provided an estimator for the variance of Sˆ JKk based on the
assumption of a multinomial distribution of ( f o , f1 , K, f n ) and application of the deltatechnique (Kendall and Stuart 1969).
The CHAO1 and CHAO2 estimators
Chao (1981) and later Chao and Lee (1992) proposed the following estimator
9
(3)
S
f
SˆCHAO1 = OBS + 1 × γ%2
Cˆ1 Cˆ1
where Cˆ1 = 1 − f1 / ncatch is an estimate of the missed species, and γ% is an approximation
to the coefficient of variation of fj , (j = 1,...., n). Chao and Lee (1992) also suggested the
alternative in (4) as an improvement over (3)
(4)
S
f )
SˆCHAO 2 = obs + 1 × γ 2
Cˆ 2 Cˆ 2
−1 −1
)
where Cˆ 2 = 1 − f1 − 2 f 2 × ( ncatch − 1) ncatch
, and γ is a potentially better approximation to
γ than γ%.
SˆCHAO1 and SˆCHAO2 are both non-linear functions of f1 ,K , f n . Hence an estimator
of their sampling variances can be obtained by applying the same assumptions and
asymptotic approximations as for the jackknifed variance estimators. Details are in Chao
and Lee (1992).
The CHAO3 estimator
Chao (1989) - in recognizing that f1 and f 2 are the two most influential
incidence frequencies for estimating richness - proposed the following estimator which
we have modified slightly to make it robust against zero-valued frequencies
SˆCHAO 3 = SOBS
(5)
(f )
+
* 2
1
2 f 2*
f1* = ( SOBS − f1 − f 2 ) ×
f1
f1 + f 2
⎛
f1 ⎞
f 2* = ( SOBS − f1 − f 2 ) × ⎜1 −
⎟
f1 + f 2 ⎠
⎝
10
A variance estimator is obtained by the same procedure as for the generalized jackknifed
estimator.
The Beta-binomial estimator (BBIN)
The observed incidence vector f can be viewed as a zero-censored outcome of
draws from a beta-binomial distribution where the probability of success (incidence of a
species) varies from plot to plot. Accordingly, Dorazio and Royle (2003) proposed a
maximum likelihood estimate of f 0 ≡ S − SOBS under this model. By adding this estimate
to the observed richness one obtains the BBIN estimator of richness. The estimated
variance of a BBIN estimate of richness is obtained by the aforementioned deltatechnique. Confidence intervals were derived from the profile-log-likelihood (Lloyd,
1999).
The mixed-binomial estimator (MBIN)
Mao and Colwell (2005) and Norris and Pollock (1998) viewed the vector f as a
zero-censored outcome of sampling from a mixture of k binomial distributions where k is
an unknown parameter to be estimated. As for BBIN, a maximum likelihood estimate of
f 0 ≡ S − SOBS is obtained under this model and added to the observed richness to give
the MBIN estimate of richness. Procedures for estimating the sampling variance and
confidence intervals of a MBIN estimate were similar to those for BBIN.
The gamma-mixed Poisson estimator (GPOI)
A zero-truncated gamma-mixed Poisson distribution (negative binomial) has been
suggested by several as the generating distribution for f (Chao and Bunge 2002, Chao
and Lee 1992, Efron and Thisted 1976, Fisher, Corbet, and Williams 1943). The
probability of ‘missing a species’ ( P0 ) is then estimated by method of maximum
likelihood under this model and used to inflate the observed richness by dividing it with
a factor 1 − P0 . As for the other model-based estimators, a variance estimator for a GPOI
estimate was obtained by application of the delta-technique. We imposed a restriction on
the location parameter of the implied gamma distribution in order to avoid absurdly high
estimates of richness. A robust estimation procedure proposed by Chao and Bunge (2002)
was used when either the estimated richness or the estimated variance became aberrant.
11
The mixed-Poisson estimator (MPOI)
Here the assumption is that f arises from draws from a mixture of k zero-truncated
Poisson distributions otherwise the estimation process is in principle the same as for
MBIN.
Finite population corrections
Our richness estimators either explicitly or implicitly assume an infinite area of
the sampled population of trees. To account for the finite area of the two compartments,
we reduced all estimates of richness (except the observed) in the following way:
(6)
SˆM′ = SOBS + (1 − f pc )( SˆM − SOBS )
where M stands for one of the nine richness estimators (Schreuder, Lin, and Teply 2000,
Valliant, Dorfman, and Royall, 2000) . This correction ensures that as f pc → 1 we get
Sˆ ′ → S
as required.
M
OBS
Evaluating estimator performance
A main objective of our study was to assess the performance of estimators of tree
species richness in the context of a forest inventory with fixed-area plots. Walther and
Moore (2005) provide a recent review of performance statistics. Our assessment is
derived from estimates of i) bias (estimated value minus the true value) ii) precision
(standard error of an estimate), and iii) accuracy (overall discrepancy between an
estimate and a nominal value). We use three statistics to quantify accuracy: 1) the mean
absolute difference between an estimate and the true value (Mad), 2) the proportion of
estimates within 10% of the true value (δ10 ) , and 3) the proportion of estimated 95%
confidence intervals that includes the true value ( pCI ) . The estimator with the lowest
Mad, the highest δ10 , and pCI closest to 0.95 for a given sampling design would be our
recommendation to the forest inventory community if the achieved levels of these
statistics were otherwise acceptable. No absolute threshold can be given for acceptable
but we surmise that an absolute error in excess of 25% and an estimate with a confidence
intervals having less than 70% of the nominal coverage would probably be useless as
12
input to a decision making process. We use a loss function (L) in (7) to compute a
performance score for each estimator (M) across all site specific sample designs:
(7)
L ( M | site ) =
∑
d ∈D ( site )
⎧ Mad
⎫
nd × p d ⎨
+ 100 ( (1 − δ10 ) + 0.95 − pCI ) ⎬
⎩ S
⎭
where d is a design in the set (D) of all site-specific designs, S is the true richness, pd is
the fraction of the site area sampled by design d, and the three accuracy statistics are all
design specific estimates. Several variants of the above loss function was tried (not
shown) with next to no consequence on the ranking of estimators. A simple average of
site-specific rankings in terms of the loss computed from (7) was used as a measure of
overall performance.
Plot effects
Plot size effects will depend on the spatial distribution and size of the trees of
each species, the sampling design and the spatial lay-out of a plot. In practical inventory
applications the important design issue is the expected effect of a change in plot size on
estimates of S and whether the expected effect is statistically significant when allowing
for sampling errors. We use Hotelling’s multivariate T2 test to assess the significance of
plot effects across sample size (Rencher, 1995). Specifically, we tested the null
hypotheses of no difference between the results obtained with 156 m2 and 400 m2 plots
and between 400 m2 and 625 m2 plots. A follow-up T2 test of plot × sample size
interactions was done for all significant plot effects.
RESULTS
Bias
Observed richness was, as expected, downward biased (Tables 1 and 2). For a
given fraction of the area sampled the bias was two to three times larger in WGHAT than
in BCI, a reflection of the much higher number of individual trees sampled in BCI. With
10% of the area sampled in WGHAT the bias was –56% as opposed to –20% in BCI. It
will take a sampling of about 18% of the BCI compartment to bring the bias below 15%.
13
The choice of richness estimator is clearly important (Tables 1 and 2). The best estimator
reduces the observation bias to between –1% and –15% in WGHAT and to between 1%
and –11% in BCI. Even the worst estimator produced an estimate with less bias than the
observed richness.
Estimates of bias varied by more than 40% between the best and the worst
estimator. Least performing were MPOI, BOOT, and MBIN. The only estimates with a
positive bias came from CHAO3, BBIN, and GPOI. Results for CHAO2 are not reported;
they were almost indistinguishable from CHAO1 results. The jackknifed estimates were a
mixture of first- (23%), second- (53%) and third- (~21%), and fourth-order (3%)
generalized estimates.
The ranking of several estimators in terms of absolute bias varied with sample
size. In WGHAT the BBIN estimator was best for n ≤ 15 whereas CHAO3 was best for n
≥ 20. The GPOI estimator consistently ranked second or third. The performance of
CHAO1 improved with increased sample size. In BCI the estimator with the lowest
average bias was CHAO3 for sample fractions below 3%, and CHAO1 when 3% to 5%
of the area is sampled. At higher sample fractions GPOI or JKk were best.
[TABLES 1 AND 2 HERE]
Precision
Estimates of standard error of the richness estimates are in Tables 1 and 2. For
one group of estimators (OBS, BOOT, and MPOI) the estimated error was 8% or less
across all designs and sites. Richness estimates from JKk, CHAO3, and BBIN had the
highest estimates of standard error. PET, CHAO1, MBIN, and GPOI produced site- and
sample size dependent estimates of error.
Estimates of error should, ideally, match the actual error observed in repeated
sampling. For PET the estimates of error were conservative (at least 25% too large)
whereas estimates from BOOT, JKk, and GPOI appear liberal (estimated errors are at
least 25% too small). Only error estimates from OBS and CHAO1 were within 25% of
the empirical error observed in repeated sampling. The reliability of error estimates from
CHAO3, BBIN, MBIN, and MPOI appears to depend on either site, sample size or both.
14
The proposed estimator of error for the observed richness appears attractive. In
contrast, an error estimate based on the assumption of a Poisson distribution of the
number of observed species would underestimate the empirical error by at least 70%.
Accuracy
Mean absolute differences between the estimated and the actual tree species
richness (Mad, Tables 3 and 4) mirrored, by and large, the results on (absolute) bias.
Again, the lowest values of Mad were obtained with GPOI and CHAO3 in WGHAT and
with GPOI, CHAO1, and JKk in BCI. Least performing on both sites were OBS, BOOT
and MPOI. A fluctuating performance was noted for CHAO1 in WGHAT and CHAO3 in
BCI. The performance of BBIN is best described as erratic. At comparable fractions of
the area sampled Mad in WGHAT was about two and a half to three times higher than in
BCI, a difference attributed, as before, to the difference in the number of sample trees.
[TABLES 3 AND 4 HERE]
The proportion of the richness estimates that were within 10% of the true value
varied among estimators, site, and sample design (Tables 3 and 4). If one wish to see
80% of the estimates within these limits, then no combination of estimator and design
could meet this standard in WGHAT. In BCI, however, a combination of 156 m2 plots, a
sample size greater than 120, and the CHAO1 estimator would. For GPOI, PET, and JKk
the sample size would need to be about three times larger in order to reach the same
standard.
Estimated 95% confidence intervals failed in most cases to include the true
richness (Tables 3 and 4). A coverage rate better than 75% was only reached by four
estimators (CHAO1 and CHAO3 for n ≥ 25, BBIN for n ≤ 15, and GPOI for n ≤ 20) in
WGHAT, and by three in BCI (CHAO1 for n = 140 and plot-size 156 m2, CHAO3 for n
< 100, and GPOI for n > 120).
Overall performance ranking
By combining three statistics of accuracy, and the proportion of area sampled into
a single indicator of performance we obtained the estimator ranking in Table 5. The
15
ranking of nine estimators was very similar across the two sites; the one exception was
for BBIN. GPOI was top-ranked, while second and third place was taken by CHAO1 and
JKk. Bottom ranks were given to MPOI and OBS.
[TABLE 5 HERE]
Plot effects
A larger plot contains, on average, more species than a smaller plot but the
increase will be less than suggested by the ratio of areas unless the tree species are
distributed completely at random throughout the study area. This plot-size effect modifies
the probability of finding a species in a plot, the vector of observed incidences, and
consequently, the estimate of richness. Plot effects were expected to diminish with
increasing sample size. In BCI the average number of species in a 156 m2 plot was 30
(±7). A plot of 400 m2 contained on average 50 (±15) and a 625 m2 plot 65 (±10). Hence,
the increase in the number of species was only 64% viz. 54% of the increase expected
under complete spatial randomness of species specific tree locations. Similarly, estimates
of the probability of finding a species in a 400 m2 plot was as rule less than expected
from a direct binomial scaling of the probabilities for a 156 m2 plot (Figure 5). From the
difference between the two probabilities we derived an estimate of the intra-plot species
correlation (clustering) of 0.14 for the 156 m2 plot. For 400 m2 plots this correlation was
close to zero.
[FIGURE 5 HERE]
The number of species found in just one or two plots has a disproportionate effect
on most estimates of species richness. Their numbers declined, as expected, with plotsize and sample size (Figure 6). Compared to a 400 m2 plot there was an addition of
about seven species that would have been observed only once in sampling with 156 m2
plots. The corresponding difference for 625 and 400 m2 plots declined from about four at
n = 30 to almost zero at n = 140. Trends in the number of species seen only in two plots
were somewhat similar for n > 30 but otherwise less pronounced. However, when the
numbers of species found in one or two plots are plotted against the fraction of area
sampled (Figure 6) they appear to fall on a single line which suggests that the plot effect,
if it exists, must be similar in both cases. A T 2 -test suggest that the effect is not
significant at the 5% level of significance).
16
[FIGURE 6 HERE]
Plot effects also depend on the estimator. With BBIN and GPOI the expected
reduction in bias with an increase in plot size did not materialize. A significant reduction
in bias of about 10% (P < 0.001) could be achieved with OBS, PET, BOOT, and MPOI
by increasing plot size from 156 m2 to 400 m2. A further increase to 625 m2 had only a
smaller (5%) effect (only significant for BOOT estimates). Figure 7 illustrates typical
plot effects for three of the four estimators with a significant plot effect. Yet, when
estimates of richness were plotted against the fraction of area sampled they appear to fall
on a single line which suggests that plot effects are weak or non-existing.
[FIGURE 7 HERE]
DISCUSSION
Estimating the number of tree species in a forest community is a first step towards
quantifying an important component of forest biodiversity. The statistical estimation
problem and choice of estimator remains a challenge (Walther and Moore 2005). As we
saw, estimators differ widely in their estimates and their differences depend on the forest
community and on sample design. To paraphrase Bunge and Fitzpatrick (1993): “the
problem is quite resistant to a statistical solution, essentially because no matter how many
species have been observed one cannot refute the possibility of a large number of rare
species”. Link (2003) states it this way: “… even with very large samples, the analysts
will not be able to distinguish among reasonable models of heterogeneity, even though
these yield quite distinct inferences about the number of species…”. Stark differences
between estimates from related and similar models (e.g. GPOI, MPOI, BBIN, and MBIN)
mirror this statement. Nayak (1996) and Starr (1979) call the problem ‘non-standard’ due
to the dependency of the estimate on the unknown parameter and the data. A negative
correlation between the estimate and the true value is another statistical anomaly (Starr
1979). Mao and Colwell (2005) recently demonstrated the essence of the estimation
challenge: An artificial enriching of real data sets with a few individuals representing rare
species significantly changed estimates of richness and their confidence intervals. O’Hara
(2005) echoes these observations.
Since our study sites had many rare and just a few common tree species we
cannot a priori expect to obtain very good estimates of tree species richness from a forest
inventory. Without a universally best estimator of richness the choice must be based on
17
expected performance. We adopted a somewhat conservative approach to gauge the
performance of richness estimators by combining three statistics of accuracy and sample
size into an index of loss. These statistics are suggestive only, not ultimate judgments
(Walter and Moore 2005). Our assumption that an inflated estimate of richness is harmful
to credibility (Schreuder, Williams, and Reich 1999) is, of course, arguable. Both inflated
and deflated estimates of tree species richness can lead to complacency in the
management and protection of forest tree species diversity, albeit for very different
reasons.
Palmer (1990, 1991) confirmed high rates of positive bias in JK1 (26%) and JK2
(70%) in a mixed hardwood stand in the Duke Forest in North Carolina (U.S.A.). As in
our study, the bootstrap estimator did not produce any positive bias. Hellmann and
Fowler (1999) assessed JK1, JK2, and BOOT in five different forest communities in
Michigan. Their results are in many ways parallel to ours. While JK1 and JK2
consistently outperformed BOOT in terms of bias the high variability of their estimates
and especially those of JK2 generated non-trivial rates of overestimation. Schreuder,
Williams, and Reich (1999) also assessed the three CHAO estimators. No positive bias
was reported for sample sizes of 20 to 700 in eleven populations representing two states
(Missouri and Minnesota) and an assembly of Loblolly pine plantations in the southeastern United States. In their study CHAO3 was at par with CHAO1 viz. CHAO2, in
contrast to the large differences reported here. We surmise that differences in the ratio
f1 / f 2 and its distribution are the cause for these discrepancies. In tree species rich wet
tropical forests (He and Hubbell 2003, Pascal and Pélissier 1996) the ratio f1 / f 2 is not
only much higher but also more variable than in sub-tropical and temperate forests
(Liermann, Steel, Rosing et al. 2004).
We only studied incidence-based estimators in our assessment. Estimators based
on extrapolation of either species-area curves or species accumulation curves have a long
history in applied ecology (Engen, 1978) but have increasingly been criticized for the
lack of a sample-based framework (Bunge and Fitzpatrick 1993). A species accumulation
curve derived from a conventional forest inventory will, in most cases, not lend itself to
extrapolation. Krishnamani, Kumar, and Harte (2004), however, obtained a surprisingly
realistic estimate of 893 for the number of tree species in the Western Ghats of India from
just 48 conventional inventory sample plots. A strong relationship between plot
similarities (absence/presence of a species) and inter-plot distances was exploited. Cao,
Larsen and White (2004) also used estimators based on plot dissimilarities for estimating
bird and fish species richness in two regions of the United States of America and found
18
them to perform reasonably well. Condit, Hubbell, Lafrankie et al. (1996), however,
found that the relationship between dissimilarity and distance to vary across a population.
Our study reiterated the importance of choosing a richness estimator and a sample
design (Brose, Martinez, and Williams 2003, Bunge and Fitzpatrick 1993, Colwell, Mao,
and Chang 2004, Gimaret-Carpentier, Pélissier, Pascal et al. 1998, Keating, Quinn, Ivie et
al. 1998). Estimators based on a mixture of truncated distributions of the probability of
species incidence do not seem to justify the added computational burden and complexity
of estimation (Lindsay and Roeder 1992, Mao and Colwell 2005). At least not if the sole
purpose is for an estimation of tree species richness. The promising performance of GPOI
needs corroboration by additional studies since it has not previously been used for the
purpose of estimating tree species richness, nor has it been widely used elsewhere. The
runner-up CHAO1 (Bunge and Fitzpatrick 1993, Bunge, Fitzpatrick, and Handley
1995, Keating, Quinn, Ivie et al. 1998, Lee S.-M. and Chao 1994, Walther and Martin
2001) has a solid track record which should make it the estimator of choice until further
studies corroborate our GPOI results. Petersen’s estimator might be favoured by agencies
opposed to any risk of an inflated estimate.
Forest inventories are generally conducted with fixed-area plots. Plot-size and
shape is generally determined by considerations of cost, logistics, and statistical
efficiency. It appears that for small to moderate sample sizes the sampling variation in
GPOI and CHAO1 estimates of tree species richness swamps the importance of plot-size
and shape (Schreuder, Williams, and Reich 1999, Schreuder, Lin, and Teply 2000). Plotsize effects are otherwise manifest due to a clustering of tree species (Condit, Hubbell,
Lafrankie et al. 1996), especially if a fixed number of trees are selected at each sample
location (Gimaret-Carpentier, Pélissier, Pascal et al. 1998).
CONCLUSIONS
In tree species rich forests, with many rare, a few common species, and a weak
spatial clustering of tree species at small spatial scales, a conventional forest inventory
with fixed-area plots can produce reasonable incidence-based estimates of tree species
richness when the estimator is carefully chosen. A gamma-Poisson estimator appears
most promising but until corroborated by other studies Chao’s 1981 estimator is
recommended.
19
ACKNOWLEDGEMENT
We thank the Smithsonian Tropical Research Institute in Panama for the use of the BCI
data and the French Institute of Pondicherry for the use of the WGHAT data. The work is
partially supported by the Alberta Ingenuity Fund, the NSERC (Canada). We are grateful
to R. Condit, R. Foster, S. P. Hubbell and the numerous volunteers in the Center for
Tropical Forest Science for their contributions to the BCI data and to the Karnataka
Forest Department for its support in WGHAT.
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25
TABLE 1. Relative bias in estimates of tree species richness for WGHAT. Actual and
average of estimated relative sampling errors are in parentheses (Actual/Estimated).
Table entries are in % of the true tree species richness of 93.
OBS
PET
BOOT
JKk
CHAO1
CHAO3
BBIN
MBIN
GPOI
MPOI
26
10 (3.2)
-75
(4/4)
-46
(4/11)
-69
(5/3)
-49
(19/11)
-47
(18/17)
-28
(39/71)
-2
(32/25)
-53
(16/24)
-10
(38/26)
-70
(5/6)
Sample size (% Area sampled)
15 (4.8) 20 (6.4) 25 (8.0)
-69
-64
-60
(4/4)
(4/4)
(4/4)
-38
-34
-29
(4/11)
(4/10)
(3/9)
-62
-57
-52
(4/3)
(4/3)
(3/3)
-39
-32
-25
(20/12) (22/13) (25/14)
-36
-28
-20
(18/18) (17/19) (17/20)
-21
-15
-6
(34/70) (31/64) (35/57)
10
20
34
26/20)
(24/20) (23/22)
-42
-40
-38
(13/36) (13/29) (13/21)
-13
-16
-17
(23/19) (17/16) (14/14)
-63
-57
-53
(5/6)
(5/5)
(5/4)
30 (9.6)
-56
(4/4)
-25
(3/9)
-48
(3/3)
-20
(24/14)
-12
(17/21)
-2
(32/62)
45
(23/24)
-34
(12/18)
-18
(12/12)
-50
(5/4)
TABLE 2. Relative bias of estimates of tree species richness for BCI. Actual and average
of estimated relative sampling errors are in parentheses (Actual/Estimated). Table entries
are in % of true tree species richness of 301.
Plot
area
in
m2
% Area
Sampled
OBS
PET
BOOT
JKk
CHAO1
CHAO3
BBIN
MBIN
GPOI
MPOI
40
156
1.2
60
156
1.9
80
156
2.5
100
156
3.1
120
156
3.8
140
156
4.4
40
400
3.2
60
400
4.8
80
400
6.4
100
400
8.0
120
400
9.6
140
400
11.2
40
625
5.0
60
625
7.5
80
625
10.0
100
625
12.5
120
625
15.0
140
625
17.5
-43
(2/2)
-37
(2/1)
-34
(2/1)
-31
(1/1)
-29
(1/1)
-27
(1/1)
-32
(2/2)
-27
(2/1)
-24
(2/1)
-22
(2/1)
-20
(2/1)
-19
(2/1)
-27
(2/1)
-22
(2/1)
-20
(1/1)
-18
(1/1)
-16
(1/1)
-15
(1/1)
-27
(1/3)
-23
(1/2)
-21
(1/2)
-19
(1/2)
-17
(1/2)
-16
(1/2)
-19
(1/3)
-16
(1/2)
-14
(1/2)
-13
(1/2)
-11
(1/2)
-10
(1/2)
-16
(1/2)
-13
(1/2)
-10
(1/2)
-9
(1/2)
-8
(1/1)
-7
(1/1)
-36
(3/2)
-31
(2/2)
-28
(2/1)
-26
(2/1)
-24
(2/1)
-22
(2/1)
-26
(2/2)
-22
(2/2)
-19
(2/1)
-18
(2/1)
-16
(2/1)
-15
(2/1)
-22
(2/2)
-18
(2/1)
-16
(2/1)
-14
(2/1)
-12
(1/1)
-11
(1/1)
-20
(14/7)
-17
(13/7)
-14
(16/8)
-15
(6/5)
-9
(22/9)
-7
(21/9)
-15
(10/6)
-10
(18/8)
-12
(6/5)
-11
(6/4)
-8
(13/6)
-6
(8/5)
-11
(13/7)
-10
(6/5)
-5
(13/7)
-4
(11/6)
-4
(5/5)
-3
(7/5)
-21
(6/6)
-15
(5/6)
-12
(5/6)
-10
(4/6)
-8
(4/6)
-6
(4/6)
-18
(4/4)
-14
(3/4)
-12
(4/4)
-10
(4/4)
-9
(4/3)
-7
(4/4)
-16
(3/3)
-12
(3/3)
-10
(3/3)
-8
(3/3)
-6
(2/3)
-5
(2/3)
-4
(14/10)
1
(13/9)
5
(13/9)
8
(12/9)
12
(14/10)
16
(16/10)
6
(14/9)
10
(14/9)
13
(14/9)
15
(14/9)
19
(16/9)
20
(15/11)
12
(14/9)
18
(15/10)
21
(14/11)
24
(13/12)
27
(12/11)
27
(12/12)
87
(15/15)
101
(16/16)
101
(8/14)
100
(9/15)
107
(11/13)
120
(37/34)
76
(10/13)
79
(8/10)
90
(10/12)
92
(5/13)
92
(7/7)
31
(20/2)
63
(6/10)
80
(7/12)
73
(6/10)
77
(13/8)
43
(24/3)
35
(24/2)
-31
(14/9)
-31
(6/6)
-27
(5/7)
-21
(6/13)
-21
(5/8)
-12
(8/30)
-26
(6/3)
-22
(5/4)
-17
(5/8)
-15
(6/6)
-13
(6/5)
-13
(6/9)
-20
(7/5)
-15
(6/4)
-13
(6/6)
-10
(6/6)
-9
(6/11)
-7
(6/13)
4
(19/7)
10
(18/7)
9
(15/6)
10
(12/6)
10
(10/6)
11
(9/5)
-11
(7/4)
-9
(5/4)
-7
(6/4)
-6
(6/3)
-5
(6/3)
-3
(6/3)
-13
(4/3)
-9
(4/3)
-7
(3/3)
-5
(4/3)
-3
(3/3)
-2
(3/3)
-35
(5/7)
-31
(5/6)
-28
(3/5)
-28
(2/5)
-26
(2/4)
-24
(2/5)
-26
(4/6)
-23
(3/5)
-20
(2/5)
-19
(2/5)
-18
(2/5)
-17
(2/5)
-23
(3/6)
-19
(2/6)
-16
(3/5)
-16
(2/5)
-14
(2/5)
-13
(2/5)
27
TABLE 3. Relative mean absolute error of tree species richness estimates for WGHAT.
Per cent of estimates within 10% of true value (δ10 ) and coverage rates of estimated
confidence intervals ( pCI ) are in parentheses (δ10 / pCI ) .
OBS
PET
BOOT
JKk
CHAO1
CHAO3
BBIN
MBIN
GPOI
MPOI
28
10 (3.2)
75
(0/0)
46
(2/15)
69
(0/0)
50
(4/15)
47
(4/28)
42
(8/59)
24
(28/78)
56
(2/14)
31
(23/76)
70
(0/0)
Sample size (% Area sampled)
15 (4.8) 20 (6.4) 25 (8.0)
69
64
60
(0/0)
(0/0)
(0/0)
38
34
29
(3/21)
(3/21)
(5/28)
62
57
52
(0/0)
(0/0)
(0/0)
41
36
31
(5/21)
(8/28)
(11/37)
37
29
22
(5/43)
(12/58) (21/76)
34
28
27
(12/67) (18/73) (19/79)
21
24
34
(35/79) (29/63) (15/36)
53
47
41
(2/18)
(2/11)
(2/9)
21
20
19
(28/80) (26/78) (27/72)
63
57
53
(0/0)
(0/0)
(0/0)
30 (9.6)
56
(0/0)
25
(8/34)
48
(0/0)
27
(16/46)
17
(33/86)
25
(23/85)
45
(4/13)
37
(4/10)
19
(24/70)
50
(0/0)
TABLE 4. Relative mean absolute error of tree species richness estimates for BCI (in per
cent of 301). Per cent of estimates within 10% of true value (δ10 ) and coverage rates of
estimated 95% confidence intervals ( pCI ) are in parentheses (δ10 / pCI ) .
n
% Area
Sampled
OBS
PET
BOOT
JKk
CHAO1
CHAO3
BBIN
MBIN
GPOI
MPOI
40
Plot
area
in m2
156
1.2
60
156
1.9
80
156
2.5
100
156
3.1
120
156
3.8
140
156
4.4
40
400
3.2
60
400
4.8
80
400
6.4
100
400
8.0
120
400
9.6
140
400
11.2
40
625
5.0
60
625
7.5
80
625
10.0
100
625
12.5
120
625
15.0
140
625
17.5
43
(0/0)
37
(0/0)
34
(0/0)
31
(0/0)
29
(0/0)
27
(0/0)
32
(0/0)
27
(0/0)
24
(0/0)
22
(0/0)
20
(0/0)
19
(0/0)
27
(0/0)
22
(0/0)
20
(0/0)
18
(0/0)
16
(0/0)
15
(0/0)
27
(0/0)
23
(0/0)
21
(0/0)
19
(0/0)
17
(0/0)
16
(1/0)
19
(0/0)
16
(2/0)
14
(6/0)
13
(16/0)
11
(29/1)
10
(45/1)
16
(1/0)
13
(11/0)
11
(29/0)
9
(57/0)
8
(80/2)
7
(94/4)
36
(0/0)
31
(0/0)
28
(0/0)
26
(0/0)
24
(0/0)
22
(0/0)
26
(0/0)
22
(0/0)
19
(0/0)
18
(0/0)
16
(0/0)
15
(2/0)
22
(0/0)
18
(0/0)
16
(0/0)
14
(2/0)
12
(6/0)
11
(24/0)
23
(11/19)
20
(10/16)
19
(12/20)
16
(12/18)
16
(27/27)
16
(30/41)
16
(15/24)
16
(19/27)
13
(33/33)
11
(36/34)
11
(48/50)
9
(57/51)
14
(26/35)
11
(42/40)
10
(52/48)
8
(72/62)
6
(91/72)
6
(85/63)
21
(3/12)
15
(16/32)
12
(31/47)
10
(54/66)
8
(72/74)
6
(86/90)
18
(4/3)
14
(12/3)
12
(32/19)
10
(49/20)
9
(60/38)
7
(76/50)
16
(3/1)
12
(25/4)
10
(54/14)
8
(79/33)
6
(94/52)
5
(100/67)
12
(48/76)
11
(52/84)
11
(62/84)
12
(49/88)
14
(51/74)
17
(34/66)
12
(54/81)
14
(42/72)
15
(42/64)
16
(38/62)
20
(34/48)
20
(32/46)
14
(47/73)
18
(39/57)
20
(22/55)
25
(10/42)
25
(14/32)
26
(11/30)
86
(0/0)
100
(0/0)
101
(0/0)
100
(0/0)
107
(0/0)
120
(0/0)
77
(0/0)
79
(0/0)
90
(0/0)
93
(0/0)
92
(0/0)
30
(10/0)
63
(0/0)
80
(0/0)
73
(0/0)
77
(0/0)
43
(0/0)
35
(0/0)
32
(4/4)
31
(2/0)
27
(6/6)
23
(6/6)
21
(14/12)
20
(14/12)
26
(0/2)
22
(0/0)
17
(12/12)
14
(24/24)
13
(36/36)
15
(18/18)
20
(2/2)
15
(6/6)
13
(22/22)
11
(36/38)
13
(29/29)
13
(18/28)
15
(48/60)
15
(46/56)
13
(47/60)
12
(54/58)
11
(52/57)
12
(45/50)
12
(37/32)
10
(62/34)
8
(64/46)
7
(72/49)
7
(74/57)
5
(88/64)
13
(26/16)
9
(57/32)
7
(85/44)
5
(91/60)
4
(98/82)
4
(100/84)
36
(0/0)
32
(0/0)
28
(0/0)
28
(0/0)
26
(0/0)
24
(0/1)
26
(0/0)
23
(0/0)
20
(0/0)
19
(0/0)
18
(0/0)
17
(0/0)
23
(0/0)
19
(0/0)
16
(2/2)
16
(3/3)
14
(6/6)
13
(3/4)
29
TABLE 5. Relative loss function scores (normalized to interval [0;1]) and rankings of
tree species richness estimators (1 = best, 10 = worst). See EQ 7 and text for details on
loss function.
WGHAT
30
BCI
Rel.
loss
Rank
Rel.
loss
Rank
Overall
Rank
OBS
1.00
10
0.54
9
10
PET
0.59
6
0.34
5
5
BOOT
0.95
8
0.49
7
8
JKk
0.52
5
0.13
2
4
CHAO1
0.20
3
0.14
3
2
CHAO3
0.12
2
0.25
4
3
BBIN
0.28
4
1.00
10
7
MBIN
0.78
7
0.37
6
6
GPOI
0.00
1
0.00
1
1
MPOI
0.96
9
0.50
8
9
FIGURE 1. Relative number of WGHAT tree species (S%) versus relative incidence in
plots of 100 m2.
FIGURE 2. Histogram of number of WGHAT tree species (S) in 100 m2 plots.
FIGURE 3. Relative number of BCI tree species (S%) versus relative incidence in plots
of 156 m2.
FIGURE 4. Histogram of number of BCI tree species (S) in a 156 m2 plot.
FIGURE 5. Left: Probability of finding a tree species in a 400 m2 plot (p400) plotted
against the probability of finding it in a 156 m2 plot (p156). Estimates from data are in
black and estimates obtained by a binomial scaling of results from 156 m2 plots are in
gray. Right: Corresponding estimates for the 625 m2 versus 400 m2 plots.
FIGURE 6. Frequencies of tree species found in only one plot (f1) and two plots (f2)
versus sample size viz. proportion of area sampled (%Area). Plot size: 156 m2 (black),
400 m2 (medium gray), and 625 m2 (light gray). The interval of plus/minus one standard
error of an estimate is indicated by a vertical line.
FIGURE 7. Estimates of tree species richness versus sample size (left column) and
proportion of area sampled (%Area). Plot size: 156 m2 (black), 400 m2 (medium gray),
and 625 m2 (light gray). The interval of plus/minus one standard error of an estimate is
indicated by a vertical line.
31
FIGURE 1.
32
FIGURE 2.
33
FIGURE 3.
34
FIGURE 4.
35
FIGURE 5
36
FIGURE 6.
37
FIGURE 7.
38
39
40